Individual differences in the production of referential...

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Serratrice, L and De Cat, C orcid.org/0000-0003-0044-0527 (2020) Individual differences in the production of referential expressions: The effect of language proficiency, language exposure and executive function in bilingual and monolingual children. Bilingualism: Language and Cognition, 23 (2). pp. 371-386. ISSN 1366-7289

https://doi.org/10.1017/S1366728918000962

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Referential choice in bilingual and monolingual children

Individual Differences in the Production of Referential Expressions: The Effect of Language Proficiency, Language Exposure and Executive Function in Bilingual and Monolingual Children* Ludovica Serratrice, University of Reading Cécile De Cat, University of Leeds *Acknowledgments This research was funded by a grant from the Leverhulme Trust (RPG-2012-633), which is gratefully acknowledged. Special thanks to Sanne Berends for leading on data collection and coding, and to Furzana Shah for assistance with the data collection. Many thanks to Arief Gusnanto for statistical consultancy, and to the many schools who opened their doors to our project, to the participating children for their enthusiasm and their efforts, and to their parents for filling in lengthy questionnaires. Address for correspondence: Ludovica Serratrice University of Reading School of Psychology and Clinical Language Sciences Harry Pitt Building Reading RG6 7BE [email protected]

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One hundred and seventy-two English-speaking 5- to 7-year-olds participated in a

referential communication task where we manipulated the linguistic mention and the

visual presence of a competitor alongside a target referent. Eighty-seven of the

children were additionally exposed to a language other than English (bilinguals). We

measured children’s language proficiency, verbal working memory (WM), cognitive

control skills, family SES, and relative amount of cumulative exposure and use of the

home language for the bilinguals. Children’s use of full Noun Phrases (NPs) to

identify a target referent was predicted by the visual presence of a competitor more

than by its linguistic mention. Verbal WM and proficiency predicted NP use, while

cognitive control skills predicted both the ability to use expressions signalling

discourse integration and sensitivity to the presence of a discourse competitor, but not

of a visual competitor. Bilingual children were as informative as monolingual

children once proficiency was controlled for.

Keywords: referential choice, anaphora, individual differences, cognitive control,

gradient bilingualism

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One of the core aspects of human communication revolves around the choice

of linguistic expressions for referent identification, i.e. the use of proper names (e.g.

Laura), Noun Phrases – NPs - (e.g. the girl, my sister, my sister’s car) and pronouns

(e.g. she, them, someone) to talk about entities in the world. Adults, and, to some

extent, preschool and school-age children are sensitive to a number of structural,

semantic and discourse-pragmatic constraints when it comes to producing referential

expressions in a communicative context (see Serratrice & Allen, 2015, for an

overview of the acquisition of reference).

Despite a general sensitivity to the aforementioned constraints, there are

individual differences in the extent to which both adults and children rely on

perspective-taking skills to process and produce referential expressions. Taking the

perspective of a conversational partner requires the inhibition of one’s own

perspective and the shifting to that of the addressee. Recent work on adult speakers

(Ryskin, Benjamin, Tullis & Brown-Schmidt, 2015; Wardlow, 2013), and some

emerging work in child and adolescent speakers (Nilsen & Graham, 2009; Nilsen,

Varghese, Xu & Fecica, 2015; Torregrossa, 2017; Wardlow & Heyman, 2016), has

identified executive function skills, particularly working memory (WM), and

cognitive control, i.e. the ability to resolve a conflict by inhibiting an irrelevant

response and promoting relevant information, as significant predictors of individual

variation in referential communication success. The use of a referential expression

implies a choice, for example a pronoun vs. a NP. This choice arises from the

selection between different options and, at least in some cases, it is the outcome of the

resolution of a conflict between competing alternatives. For example, if the speaker

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and the addressee have different levels of access to a target referent, their mental

representations will not entirely overlap. The onus is on the speaker to inhibit a

potentially egocentric perspective and promote an addressee-friendly perspective that

will maximise the chances of convergence between the mental representations of both

speaker and addressee. This can translate into choosing a more informative NP (e.g.

the tall girl), as opposed to a more reduced and less informative expression (e.g. she).

Because conflict monitoring and resolution depend on the inhibition of irrelevant

information, the promotion of relevant information, or both, we will adopt the term

cognitive control to include both the inhibition and the promotion aspects of the

process (Teubner-Rhodes, Mishler, Corbett, Andreu, Sanz-Torrent, Trueswell &

Novick, 2016).

WM refers to the ability to store and manipulate information, and it has been

connected to perspective-taking and referential choice in at least two ways. Firstly, it

underpins the storage and updating of the interlocutor’s perspective and the

comparison of that perspective with one’s own to check for convergence (Nilsen &

Bacso, 2017; Wardlow, 2013). Secondly, it may be implicated in the use of feedback

in the case in which one of the interlocutors explicitly signals a mismatch between

their perspective and that of their conversational partner. Higher verbal WM capacity

has been shown to correlate positively with 5- and 6-year-olds ability to use an adult’s

non- verbal feedback to produce a discourse-appropriate referential expression

(Wardlow & Heyman, 2016).

A parallel line of research has singled out bilingual speakers – both older

adults and children - as having an advantage in the same executive function skills of

cognitive control that are associated with referential choice (Bialystok & Martin,

2004; Morales, Calvo & Bialystok, 2013). Whether bilinguals genuinely have

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superior WM skills compared to monolinguals, or not, is, however, not yet clear.

Some studies report no difference between bilingual and monolingual children

(Barbosa, Jiang & Nicoladis, 2017; Bialystok, Luk, and Kwan 2005; Engel de Abreu,

2011), others report an advantage for bilingual children (Morales, Calvo, & Bialystok,

2013).

In the present study we combine these two independent lines of inquiry to

investigate how degrees of exposure to/and use of English and another home

language, language proficiency in English, and executive function skills (cognitive

control and verbal WM), predict the choice of linguistic expressions in a referential

communication task in monolingual and bilingual children between the ages of 5 and

7. In the task we manipulated a linguistic factor (the discourse mention of a

competitor to the target referent) and a non-linguistic factor (the visual presence of a

competitor to the target referent) to provide new evidence on the sources of contextual

information used by children in reference production. Previous work has focused on

children’s use of deictic expressions in referential communication tasks (e.g. Nilsen &

Graham, 2009), while we were specifically interested in children’s use of anaphoric

expressions to refer to a previously mentioned antecedent.

Research including bilingual children has sometimes neglected to take into

account the SES profile of participants. This is an important limitation as SES is

known to be predictive of both language and of cognitive skills. In the present study

we therefore included a measure of SES in our analyses.

��

Adult speakers are sensitive to a number of structural and discourse-pragmatic

constraints in their referential choices. They tend to use more pronouns for referents

that are in subject position (Arnold, 2001) and/or in sentence-initial position

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(Järvikivi, van Gompel, Hyönä & Bertram, 2005), or for referents that are topics

(Anderson, Garrod & Sanford, 1983). Conversely, competent speakers tend to use

more informative referential expressions (e.g. proper names and indefinite NPs) when

the referent is new to the discourse (Gordon, Hendrick, Ledoux & Yang, 1999), or

when the use of a pronoun might lead to potential ambiguity (Arnold, 2008). Adult

speakers generally can take the perspective of their listener into account, and they

choose their referential expressions accordingly. Perspective-taking is predicated

upon the ability to distinguish between what is in the common ground (Clark, 1992),

and therefore shared knowledge between speaker and listener, and what is in the

privileged ground, i.e. knowledge that is only accessible to the speaker. The common

ground can either be established perceptually, i.e. when it includes referents that are

visually accessible to both interlocutors, and/or it can be established linguistically via

the use of discourse-appropriate referential expressions.

Competent adult speakers typically engage in modelling their addressee’s

perspective to produce a referential expression that is optimal for their conversational

partner (Hendriks, Englert, Wubs & Hoeks, 2008). In essence the assumption is that

competent speakers maintain their onw mental representation of their addressee’s

mental representation. However, the extent to which these meta-representations

always require an effortful and intentional commitment on the part of the speaker, and

whether they necessarily rely on explicit Theory of Mind skills, is debated in the

literature (Horton & Brennan, 2016).

Even before they have a fully developed Theory of Mind, three-year-olds are

already at least partly sensitive to the same constraints that regulate referential choice

in adult speakers (see Allen, Hughes & Skarabela, 2015, for a review). Pre-school

children are more likely to omit arguments, or use reduced expressions, when they are

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part of the common ground either through joint attention (Skarabela, 2007), previous

linguistic mention (Allen & Schröder, 2003; Clancy, 2003; Guerriero, Oshima-

Takane & Kuriyama, 2006; Stephens, 2015), or prior mention and/or perceptual

availability (Campbell, Brooks & Tomasello, 2000; De Cat, 2011; Matthews, Lieven,

Theakston & Tomasello, 2006; Rozendaal & Baker, 2010; Salazar Orvig et al., 2010a;

Salazar Orvig et al., 2010b).

At the same time, children are notoriously less capable than adults when it

comes to taking their listener’s perspective into account and to adjusting their

referential choices accordingly. This has been observed in production studies in pre-

schoolers (De Cat, 2011, 2015), in five-year-olds (Theakston, 2012), and in six-years-

old (Serratrice, 2008) when children need to provide a referential expression, and up

to adolescence in comprehension where participants need to make a choice between

potential referents (Dumontheil, Apperly, & Blakemore, 2010).

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It is becoming increasingly apparent that there are individual differences in the

degree of perspective-taking abilities, and that this variation may correlate with the

ability to interpret referential expressions in discourse-pragmatic appropriate ways

(Brown-Schmidt, 2009; Lin, Keysar & Epley, 2010; Ryskin et al., 2015). Studies on

adults have focused on the relationship between perspective-taking abilities - indexed

by referential choice - and cognitive control and WM - two core components of

executive function. There is some additional evidence that cognitive control also

plays a role in perspective-taking and referential interpretation in pre-school children.

In two referential communication studies with three- and five-year-olds, Nilsen and

Graham (2009) reported that performance on a cognitive control task significantly

predicted comprehension accuracy for both the younger and the older children.

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However neither WM nor cognitive control were predictive of accuracy in a

production task in which the five-year-olds had to provide a disambiguating adjective

to identify a referent in the privileged ground condition. Nilsen and Graham (2009)

speculated that this non-significant finding could be due to the fact that their measure

for assessing children’s perspective taking (i.e. the number of adjectives in the

common ground condition) was not sufficiently sensitive to reveal the impact of

cognitive control.

Some of the adult studies point to a positive correlation between cognitive

control skills and perspective-taking abilities in the online interpretation (Brown-

Schmidt, 2009; Lin et al., 2010) and production of referential expressions (Wardlow,

2013), but others have failed to replicate this finding with monolingual and bilingual

adults in a spatial perspective-taking task (Ryskin, Brown-Schmidt Canseco-

Gonzalez, Yiu, & Nguyen, 2014), and with children with ADHD in a referential

communication task (Nilsen, Mangal & Macdonald, 2013).

Verbal WM (WM) has also been recently linked to individual differences in

perspective-taking skills in the production of referential expressions in monolingual

adults (Wardlow, 2013). Referential choice requires the speaker to focus on those

conceptual features that make the target different from potential competitors that may

or may not be accessible to the addressee. This evaluation process relies on the

storage in memory of the features of the target and it additionally requires a

comparison with the features of the competitors. This is a complex set of operations

that involve both the storage and the manipulation of information. In essence these

demands are comparable to those of a WM task where the information must be

retained in memory while being subjected to additional operations. Adopting a

computational modelling approach, Hendriks (2016) has argued for individual

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differences in WM capacity and processing speed as predictors of informativity in

referential choice. Hendriks (2016) reports on a series of computational simulations

where the manipulation of WM capacity in the network led to significant differences

in the use of pronouns vs. NPs to refer back to a potentially ambiguous antecedent

(van Rij, 2012). In the low WM model there was a significantly higher proportion of

underspecified and underinformative pronouns than in the high WM model where

more pragmatically adequate NPs were used.

The role of verbal WM has not yet been explored in connection with

referential choice in bilingual children. In monolingual children, Nilsen and Graham

(2009) did not find WM to be predictive, possibly because of the relatively low task

demands, but Wardlow and Heyman (2016) found it to be positively correlated with

5- and 6-year-olds’ ability to benefit from adult non-verbal feedback in a referential

production task. Children with higher WM improved their use of discourse-

appropriate referential expressions in the course of the experiment when they received

feedback that they were being uninformative. In a sample of monolingual German-

speaking 8- to 10-year-olds Torregrossa (2017) also found a positive correlation

between WM - indexed by backward-digit-span scores - and the discourse-

appropriate use of demonstrative pronouns in a story-telling task pronouns. In the

light of Wardlow’s (2013) preliminary findings with adult speakers, Torregrossa’s

(2017) findings with 8- to 10-year-olds, and the results in the feedback condition for

the 5- and 6-year-olds in Wardlow and Heyman’s (2016) study, it is theoretically

interesting to test whether the relationship between choice of referring expressions

and verbal WM generalizes to bilingual child speakers

��

A parallel but independent line of research has shown, albeit not

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uncontroversially (see Valian, 2015), that cognitive control is one area in which

bilinguals may have an advantage over monolinguals (Bialystok, 2015). If bilingual

children do have an advantage when it comes to inhibiting information that is in their

privileged ground and promoting information in the common ground, and if this kind

of cognitive control is conducive to referential communication, it follows that

bilingual children should, in principle, be more successful in choosing discourse-

appropriate linguistic expressions in a referential communication task that requires

cognitive control. To date no studies have directly investigated whether individual

differences in cognitive control and WM confer an advantage to young bilinguals

when it comes specifically to referential choice. The literature on referential

expressions in bilingual children and adults has principally focused on the issue of

cross-linguistic influence, and on whether the interpretation of third person pronouns

is affected in a null-subject language when the other language has obligatory overt

subjects (Serratrice & Hervé, 2015). More recently some studies with infants and

young children have reported a bilingual advantage for sensitivity to referential cues

(Fan, Liberman, Keysar, & Kinzler, 2015; Liberman, Woodward, Keysar, & Kinzler,

2017)

Although superior cognitive control skills may put bilingual children in a

privileged position in terms of perspective-taking and referential choice, other factors

must also be considered as predictors of discourse-appropriate linguistic choices. The

bilingual language experience is, by its very nature, distributed across language, and –

at least in relative terms - bilingual children receive proportionally less input in each

language that monolingual children. Although relative amount of exposure is only an

indirect and imperfect approximation of input quantity (Carroll, 2017; De Houwer,

2014; Hurtado, Grüter, Marchman & Fernald, 2014), it has repeatedly been shown to

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correlate robustly with measures of language proficiency (Hoff, Welsh, Place &

Ribot, 2014; Unsworth, 2013).

It is plausible to expect a positive correlation between overall language skills

and the ability to select discourse-appropriate referring expressions. Hence, whatever

advantage superior cognitive control skills might confer to bilinguals when it comes

to referential choice – if any – it may be offset by lower language proficiency when

compared to monolingual children. Ryskin et al. (2014) make a similar claim to

account for the lack of a bilingual advantage in a spatial perspective-taking task with

adults. Some evidence that language proficiency may play a role comes from a

referential communication study (Fan, Liberman, Keysar, & Kinzler, 2015) which

also included measures of language proficiency (receptive vocabulary), cognitive

control, and fluid intelligence, in a group of monolingual 5-year-olds and two groups

of age-matched children who were either bilingual, or exposed to a multilingual

environment. The only significant effect was that of group with both the bilingual and

multilingual exposure children outperforming the monolinguals. Crucially the three

groups did not differ in terms of receptive vocabulary, and therefore it remains to be

seen whether bilinguals with lower language skills than monolinguals might be

adversely affected in a linguistic task.

Another variable that may potentially affect children’s linguistic and cognitive

performance is SES. SES is a complex construct and it is considered a proxy for

access to a range of economic, educational and occupational resources (Hauser &

Warren, 1997; McLoyd, 1998). Although there is a vast and expanding literature on

the relationship between SES and language and cognitive development, attributing a

causal role to SES in child development is not straightforward because SES is a

multifaceted notion and so are language and cognition (Duncan & Magnuson, 2012).

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For example, SES has been shown to affect vocabulary size but not utterance length

(Hoff-Ginsberg, 1998), grammar but not pragmatic development (Wells, 1986), and

the effects are greater for expressive than receptive vocabulary (Snow, 1999).

In monolinguals the complex relationship between linguistic and cognitive

development and SES is well documented (Hackman & Farah, 2009; Hackman,

Gallop, Evans & Farah, 2015). When it comes to bilingual children, there is inevitably

an added layer of complexity. In bilingual populations SES also has a predictive role

on language and cognitive skills, although it is not often easy to tease apart the

relative contribution of bilingualism and SES. In many studies there are significant

cultural differences between the bilingual and the monolingual groups, and the

immigrant status of the bilinguals may present an additional confound. A number of

studies have recently tried to disentangle SES from bilingualism (Calvo & Bialystok,

2014; Carlson & Meltzoff, 2008) and the main finding seems to be that both

bilingualism and SES independently account for the variance observed in linguistic

and cognitive tasks. The relationship between SES, bilingualism, and language and

cognitive performance is however complex (Gathercole, Kennedy & Thomas, 2015)

and is mediated by language exposure, age and the specific aspect of language (e.g.

vocabulary vs. grammar), or of non-verbal cognition being tested.

��

To date, the relationship between perspective-taking skills, cognitive control,

verbal WM, and referential choice has mostly been studied in the context of online

comprehension. Studies investigating the predictive role of executive function skills

in production have reported mixed results (Nilsen & Graham, 2009; Wardlow Lane,

2013; Ryskin et al., 2015; Torregrossa, 2017; Wardlow & Heyman, 2016).

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The first aim of the present study is to test whether cognitive control, as

measured by the Simon task, and verbal WM, as measured by backward digit recall,

are predictive of referential choice in a production task in which child participants

need to build a complex situation model and identify a target referent in settings in

which we manipulate the presence of discourse and visual competitors. The prediction

is that the Simon task score and the backward digit recall score will correlate

positively with the informativeness of the participants’ referential choices.

The second aim of the present study is to investigate the contribution of language

experience to perspective-taking abilities and referential choice. English-speaking

monolingual children and bilingual children with varying degrees of exposure to a

language other than English (henceforth the home language) are therefore included in

the study. Language experience is conceptualized here both in terms of cumulative

amount of exposure and use of the home language (Bilingual Profile Index, BPI, De

Cat, Gusnanto & Serratrice, 2017; De Cat & Serratrice, under review), and in terms of

language proficiency as measured by the Articles sub-test of the Diagnostic

Evaluation of Language Variation (Seymour, Roeper & de Villiers, 2003), a dialect-

neutral assessment for 4- to 9-year-olds, that minimizes the effects of language

exposure differences in bilingual and bicultural children. We expect that children with

better language proficiency – which is in turn likely to be predicted by the amount of

exposure and use of English – will be more sensitive to the presence of discourse and

visual competitors. It is also conceivable that language experience and language

proficiency would interact, such that bilingual children might display an advantage

only if their English proficiency falls within the range of their monolingual

counterparts – as shown by Fan et al. (2015).

Finally, studies of perspective-taking skills have typically investigated the

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comprehension and use of NPs containing disambiguating size or colour adjectives

(e.g. the small duck, the red square) that directly pick out an entity in a visual display

and are therefore not anaphoric (e.g. Nilsen & Graham, 2009; Wardlow & Heyman,

2016). In contrast, in the present study we are focusing on the use of anaphoric

expressions, i.e. third person pronouns vs. NPs, and on how the discourse and visual

contexts determine the choice of a referential expression for a target referent in the

presence of one or two antecedents that may be either visually present, linguistically

mentioned, both, or neither.

The experiment is modelled on the studies in Fukumura, van Gompel and

Pickering (2010) with monolingual adult participants where they manipulated the

linguistic mention and the visual presence of a competitor to a target referent.

Although Fukumura et al. (2010) did not address this issue, the use of an NP in

conditions in which a pronoun is ambiguous should – at least partly – be predicted by

cognitive control and verbal WM. Those participants that are more successful at

inhibiting their egocentric perspective, and have better WM resources to deal with a

complex scene, should be those that are sensitive to the presence of a discourse and

visual referent that is in competition with the target.

Our prediction is that, if - similarly to adults – children are sensitive to both

the linguistic and the non-linguistic features of the context in creating a discourse

model, they will produce more informative referential expressions, i.e. full NPs (e.g.

the princess, the cowboy) when the competitor is previously mentioned and when it is

visually present.

SES will be included as a predictor in the analyses alongside measures of

language proficiency, language exposure and use, cognitive control and verbal WM,

to assess the contribution that these child-internal factors might make to the use of

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anaphoric expressions in a demanding language production task.

��

!��

After receiving ethical approval for the study by the University Research Ethics

Committee of the second author’s institution, children were recruited in state primary

schools in the North of England. The final sample included 172 children attending

year 1 or year 2 of primary school (between the ages of 5 and 7), all of whom were

schooled exclusively in English. Half of the children (N = 87) were also exposed to a

language other than English at home; these children will be referred to as bilinguals.

In this study we adopted a broad definition of bilingualism that reflects the typical

situation of many classrooms in the UK where children are classified as learners of

English as an Additional Language (EAL) if ‘a first language, where it is other than

English, is recorded where a child was exposed to the language during early

development and continues to be exposed to this language in the home or in the

community.’ (DfE School Census Guide 2016-2017, p.63). Because of this

inclusionary criterion, the children in our bilingual group had a wide range of

exposure (as low as 9%) to 28 different home languages: Punjabi (21% of bilingual

participants), Urdu (17%), Arabic (9%), French (8%), Spanish (6%), Bengali,

Cantonese, Catalan, Dutch, Farsi, Greek, Hindi, Italian, Kurdish, Mandarin, Marathi,

Mirpuri, Nepalese, Pashto, Polish, Portuguese, Shona, Somali, Swedish, Tamil,

Telugu, Thai, Tigrinya (languages with no percentage indicator accounted for less

than 5% of the sample). Our bilingual group was therefore deliberately heterogeneous

to capture the variability of children who are currently considered as bilingual (EAL

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learners) in multilingual classrooms in the UK, and to capitalise on the notion of

bilingualism as a continuous measure.

��

In addition to the main referential communication task that is the object of this

study, we collected information on the children’s SES, on their exposure and use of

English and the home language, and tested their proficiency in English, their verbal

WM and their cognitive control skills.

Socio�economic Status (SES). The children came from schools in a range of

different catchment areas to ensure variation in SES. We collected information on

parental education and occupation via questionnaires. Children were allocated an SES

score on the basis of the highest level of occupation or education in the household

(either mother or father). Education was coded on a five-point scale (none, primary,

secondary, further, university), and the occupational data was coded according to the

reduced method of the UK National Statistics socio-economic classification. We used

the reversed occupational data scores to make the interpretation of the association

with the educational level data more transparent, so that a higher value represents an

advantage. As expected there was a strong association between the two measures (Χ2

(4, N = 174) = 83.57, p < 0.0001). We also found a weak but significant negative

correlation between level of bilingualism as measured by the children’s cumulative

amount of exposure and use measured by the Bilingual Profile Index - as described

below- and SES as measured by parental occupation (r = −.25, p = 0.0009).

Language exposure and use. We used a parental questionnaire to estimate the

bilingual children’s relative amount of exposure and use of English and of the home

language. The questionnaire, which includes both current and cumulative estimates of

the amount of exposure and use, is modelled on the BiLEC (Unsworth, 2013). The

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parents (usually the mother) completed the questionnaire in English, Bengali, Punjabi

or Urdu with the help of a bilingual assistant. They were asked to quantify the amount

of their child’s current exposure and use of the two languages on a typical school day,

at weekends, and during holiday periods. School days were divided into slots of one

hour before and after school during which children were exposed predominantly to

English. It is possible that children may have used the home language with some

same-language peers at school but because parents – and not teachers – were asked to

complete the questionnaire, we did not have access to this information and we

conservatively assumed that during school hours children only heard and used

English. Parents were asked about all of the child’s interlocutors, and to estimate on a

five-point scale how often they addressed the child in the home language (never,

rarely, half of the time, usually, always). We later converted the scores into discrete

percentage bands ranging from 0 (never) to 100% (always). Parents were also asked

to recall age of first exposure to English. To calculate the current relative amount of

exposure to English and the home language for a given child we extrapolated the

number of hours that the child spends with each interlocutor on a yearly basis, and we

multiplied this figure for the percentage of time the child used either English or the

home language with each interlocutor. The percentages for each of the child’s

interlocutors were added and then divided by the total number of hours of interaction

pooled for all interlocutors, if several interlocutors were present at the same time, the

estimate was divided by the number of interlocutors for the relevant time window.

The resulting was a percentage expressing the relative amount of input for English

and the home language. We applied the same method to the calculation of a relative

measure of child’s output, i.e. use of English or the home language. For the

cumulative amount of input/output in each language we firstly calculated the number

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of months of home language use only, i.e. before children were exposed to English –

this was 0 for the simultaneous bilingual children – we then multiplied the number of

months of bilingual exposure by the proportion of current input/output. The resulting

figure is the total number of months equivalent to full-time exposure to the home

language.

The use of parental questionnaires to collect information on quantity and

quality of child-directed input has obvious limitations and has lately come under

critical scrutiny (Carroll, 2017). Although we acknowledge the constraints of this data

collection method, we are also confident that it is a pragmatic solution whose validity

and robustness have been repeatedly confirmed (De Houwer, 2017; Paradis, 2017).

Current and cumulative measures of input and output in the home language

were highly correlated in our sample (current input and output: r = .90, p < 0.0001;

cumulative input and output: r = .95, p < 0.0001). Because we wanted to use both

dimensions of the language experience as predictors in our analysis but needed to

avoid collinearity for modelling purposes, we used Principal Component Analysis

(PCA) to decorrelate the two measures and create a composite score of cumulative

input and output which we call the Bilingual Profile Index (BPI, De Cat et al., 2017;

De Cat & Serratrice, under review). The PCA of cumulative input and cumulative

output yielded two principal components, the first of which captured 98% of the

variability (given the strength of the correlation between the two cumulative

measures). The BPI scores correspond to the loadings of that first component,

reversed (so that a higher score corresponds to more experience in the home

language) and aligned with a score of 0 for monolinguals. The BPI can be interpreted

as a cumulative and gradient measure of a bilingual child’s experience of their home

language, effectively close to the number of full-time months of exposure corrected

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for any imbalance between exposure and use. The range of the BPI in our sample is

from 0 to 96.

Language proficiency. We used the Articles sub-test of the Diagnostic

Evaluation of Language Variation - DELV (Seymour et al., 2003) as a measure of

language proficiency in English, the language of schooling. The DELV is a language

assessment of syntax, semantics, pragmatics and phonology for children between the

ages of 4 and 9. This test was specifically developed to neutralize dialectal differences

and it focuses on language structures that are common to all children from English-

speaking backgrounds regardless of the particular variety of English they speak. We

chose the Articles sub-test as an independent measure of language proficiency as it

taps into some of the same discourse-pragmatic skills that are required for the

appropriate use of referential expressions.1

Verbal working memory (WM). We used the Backward Digit Span task from

the Wechsler Intelligence Scales for Children (Wechsler, 1991) as a proxy measure

for children’s verbal WM capacity. The backward digit span was administered

according to the WISC-IIIUK instructions: for each digit span the experimenter

administered two trials, regardless of whether the first trial was passed or failed, and

discontinued the test after failure on both trials of any item. Backward digit recall is

one of three complex memory span measures (the other two being listening recall and

counting recall) that in a confirmatory analysis were shown to load onto one single

1 Performance in this proficiency task is significantly correlated with performance on

other language proficiency measures collected as part of our larger study including

the School-Age Sentence Imitation Task (Marinis, Chiat, Armon-Lotem, Gibbons &

Gipps, 2010). See De Cat & Serratrice (under review, https://osf.io/wkgv7/) for

details.

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factor by Gathercole, Pickering, Ambridge and Wearing (2004). Unlike forward digit

recall, which only requires the storage and immediate recall of a sequence of spoken

items and taps into the phonological loop, backward digit recall implies both the

phonological loop, for the storage of items, and the central executive, for the

additional processing in the reversing of the digits.

Cognitive control. Children were administered a computer-based version of

the Simon task (Simon & Wolf, 1963) programmed and run via E-Prime. The Simon

task is considered a complex response inhibition task (Garon, Bryson & Smith, 2008).

because it involves moderate WM demands in addition to the inhibition of a prepotent

response. Participants need to hold a rule in mind (press the left button when you see

x, press the right button when you see y), respond according to this rule (physically

press the key), inhibit a prepotent response when the rule changes and respond

accordingly (press left button when you see y, press the right button when you see x).

The Simon task is one of many complex inhibition tasks that have been used

in the developmental literature to measure children’s ability to inhibit a prepotent

response while responding to a salient conflicting response option (see Garon et al.,

2008 for a comprehensive review). With specific reference to the bilingual-

monolingual comparison, previous studies have shown that bilingual children

outperform monolingual peers only in tasks that assess the interference suppression

component of cognitive control (Bialystok & Shapero, 2005; Qu, Low, Zhang, Li &

Zelazo, 2016), but not in tasks that assess response inhibition alone (Martin-Rhee &

Bialystok, 2008).

Children sat in front of a 15.6” computer screen and used an E-Prime serial

response button box with colour-coded buttons (red on the left and green on the right).

Children started with 8 practice trials followed by 48 test trials; there was no neutral

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condition in which the coloured square would appear in the middle of the screen.

Accuracy and Reaction Times (RTs) were automatically recorded by E-Prime. The

index of cognitive control abilities used as a predictor in the present study

corresponds to the modelled score in the Simon task, i.e. children’s score adjusted for

age, SES, bilingual experience (indexed by the BPI), and accuracy at the previous

trial.2 These correspond to the significant predictors of a Cox Proportional Hazard

regression analysis, as reported in detail in De Cat et al. (2017). The Cox PH model

captures response accuracy and speed within the same analysis, so the resulting score

combines both aspects of children’s performance.

Table 1 provide descriptive statistics for the monolingual and bilingual groups:

Insert Table 1 here

��"��

Following the design of the studies in Fukumura et al. (2010), the experiment

manipulated the visual presence and the linguistic mention of a competitor to a target

referent in a 2x2 design in four conditions: competitor present and mentioned,

competitor present and not mentioned, competitor absent and mentioned, competitor

absent and not mentioned. There were five items in each of the four conditions and

ten filler items. Each experimental item consisted of a set of two coloured

photographs of iconic Playmobil characters (e.g. fireman, cowboy, ghost, queen),

while the fillers included coloured geometric shapes and animals. Both the first and

the second photograph in the experimental set always included the target referent (e.g.

a fireman). In the competitor present conditions another referent of the same gender

2 The modelled score was obtained using the predict function of the survival package

in R (version 2.38.3), which was used for the analysis.

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also appeared in both photographs (e.g. a fireman and a pirate). Half of experimental

items contained characters of feminine gender, and the position of the target and the

competitor was counterbalanced throughout the experiment.

See Figures 1 and 2 for examples of experimental items in the competitor

visually present or absent conditions, and the Appendix for a full set of experimental

and filler items.

Insert Figure 1 and Figure 2 here

The first photograph in each set was presented alongside a digitally recorded

sentence spoken by a female native speaker of Northern British English. The sentence

was a passive whose subject contained a genitive phrase where the possessor was the

animate target referent and the possessum was an inanimate entity (e.g. The fireman’s

bed has been made). In the conditions in which the competitor was mentioned it

appeared in the passive’s by-phrase (e.g. The fireman’s bed has been made by a

pirate).

The rationale for embedding the target referent as the possessor in a genitive

phrase (e.g. The fireman in The fireman’s bed) was to reduce its accessibility and thus

generally decrease the likelihood that participants would only ever use pronouns in

their continuation. It also allowed us to tease apart sentence-initial position from

topichood. Like Fukumura et al. (2010) we also wanted to ensure that the bias

towards using a pronoun for a highly salient subject antecedent would not completely

obliterate the role of the visual context. The photographs were embedded in a

PowerPoint presentation. The second picture appeared after the first had disappeared

off the screen and was accompanied by the pre-recorded prompt “And now…“.

!��

The children were tested on school premises. Two female experimenters took

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part in the task; experimenter A sat next to the participant; the participant sat in front

of a laptop computer and the two were separated by a divider so they could not see

what the other was looking at but they could see each other. Experimenter B

introduced the task to the participant as a communication game and explained that the

aim was to give instructions to experimenter A so that she could re-create the scenes

in the child’s pictures with the toys that she was given by experimenter B.

Experimenter B pressed the space bar on the child’s laptop on each trial to start the

experiment and to move on to the next item. Before the experiment started there were

two practice trials with feedback. No children had to be discarded for not

understanding the task. At the start of each trial experimenter B pressed the space bar

and the first picture appeared on the computer screen accompanied by the pre-

recorded linguistic description (e.g.. “The fireman’s bucket has been filled (by a

musician)”) lasting an average of 4000 ms. The space bar was pressed again at the

end of the sentence and the target picture would appear accompanied by the prompt

“And now…”. This was the participant’s cue to start giving directions to experimenter

A to arrange the toys to recreate the scene that the child would describe (e.g. And now

the fireman/he/the man is carrying the bucket). Experimenter A had the same toys

that were present in the child’s picture. When the participant had completed their

instruction they looked round the divider to see whether the experimenter’s toy

arrangement matched the photograph on their computer screen. The experimenter

remained in their seat, they showed the participant their toys and asked “Like that?”.

Whenever the participant used an under-informative pronoun, experimenter A always

chose the competitor to give the participant indirect feedback about their level of

underinformativity.

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Participants’ instructions to experimenter A were digitally recorded and

transcribed using CHAT for CHILDES (MacWhinney, 2000); utterances were later

imported into Excel and coded for the following features: mention of target referent

(1= target referent; 0 = competitor); label used (repeated name from the preamble

sentence, e.g. the king; an alternative label in the same semantic field– e.g. the prince

instead of the king; an alternative label that only matched the referent in gender, e.g.

the man instead of the king, the lady instead of the dentist); discourse integration (1=

pronouns and definite NPs anaphorically referring to the target referent- e.g.

he/she/the queen; 0 = indefinite pronouns – e.g. somebody – and indefinite NPs – e.g.

a man - that do not make clear anaphoric reference to the target).

The “discourse integration” coding operates a binary distinction between

anaphoric and non-anaphoric expressions; the “label used” coding provides a more

fine-grained distinction within different types of anaphoric referential expressions.

While the king, the prince, the man are all definite NPs, they vary along a continuum

of disambiguating information. We deliberately chose stereotypical and easily

identifiable referents for the experimental items (i.e. king, fireman, astronaut, queen,

nurse, etc.). To be maximally informative in the task, participants should ideally have

used the label that was provided in the preamble description associated with the first

photograph in the experimental pair. Using a different and less informative label

might lead to potential ambiguity that would, in turn, increase as a function of the

label’s lack of informativeness. So, in the case of a label in the same semantic field

(e.g. prince instead of king) the likelihood of ambiguity would not be as high as in the

case of a highly underspecified definite NP like the man that would give experimenter

A only a vague cue to select the appropriate target toy to reconstruct the scene, and

would be just as underinformative as a third person or an indefinite pronoun.

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�

��

Table 2 provides descriptive statistics for the results of the DELV Articles

sub-test (language proficiency), the backward digit recall task (verbal WM) and the

Simon task (cognitive control) for the monolingual and the bilingual groups. Note that

the scales are different for the three measures. For the DELV, it is accuracy

proportion from 0 to 1; for the backward digit recall it is the number of accurately

recalled digits from 0 to 4 (as a score), and for the Simon task it is an index of

cognitive control adjusted for age, SES, bilingual experience and accuracy at the

previous trial; negative scores indicate better cognitive control skills.

Insert Table 2 here

A linear regression model fitted using the lme4 package (version 1.1.11) in R

(version 3.2.4) to the overall score in the DELV Articles sub-test showed that

performance was negatively correlated with the BPI (t(168) = -2.90; p = 0.004); as

expected, bilingual children performed more poorly than monolinguals overall,

greater exposure and use of the home language was correlated with lower proficiency

scores. There was no significant effect of the BPI in the verbal WM task (t(181) = -

0.29; p = 0.77). For the Simon task the results of a Cox-P Regression model showed

a near-significant effect of group (X2(1) = 3.8, p = 0.05) and a significant effect of

home language experience over and above the effect of group, as the BPI was a

positive predictor (X2(1) = 12.13, p = 0.0005). There was however no significant

interaction between bilingualism and cue congruency, and hence no Simon effect in

the strict sense (in line with previous studies).

We conducted three analyses to address the role of cognitive control, verbal

WM, cumulative home language exposure and use, SES, and language proficiency on

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the children’s use of referential expressions. In the first analysis, following Fukumura

et al. (2010), our DV only included exact repetitions of the target referent named in

the context sentence vs. the use of third person pronouns. Two further analyses were

necessary to capture the broader picture. In the second analysis, we included all

referential expressions that made anaphoric reference to the target and investigated

their informativeness by creating a binary DV: (1) underinformative expressions:

third person singular pronouns (e.g. he/she) and underinformative definite NPs – e.g.

the man instead of the king, the lady instead of the queen; and (2) definite NPs that

were either exact repetition of the definite NP in the preamble sentence, or

semantically related labels (e.g. the prince instead of the king, the singer instead of

the musician).

The third analysis identifies the factors that predict lack of discourse

integration. We used a two-way distinction between indefinites signalling a lack of

anaphoric discourse integration (i.e. indefinite NPs and indefinite pronouns), and

pronouns and definite NPs that made anaphoric reference to the target.

We fitted generalized linear mixed models using the lme4 package (version

1.1.15) in R (version 3.4.4). The models were fitted incrementally by adding

predictors one by one and retaining them only if they improved the model fit, yielding

a significant reduction in AIC and a significant R-squared value, with model

comparison estimated by likelihood ratio tests (Baayen, 2008). In each of the three

analyses we treated item as a random factor, participant was not included as random

factor because it would compete with the fixed factors capturing participant-related

variables such as the BPI, SES or proficiency. We tested for the significance of the

following fixed factors: the presence/absence of a discourse or a visual competitor,

the Simon task score (cognitive control), the backward digit recall score adjusted for

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age and proficiency (verbal WM), the DELV Articles sub-test score (language

proficiency), the BPI score (cumulative home language use and exposure), the SES

score, and age (in months). Age and Simon task scores were centered to facilitate the

interpretation of the models. The following interactions were also tested in all

analyses: visual competitor x discourse competitor (yielding the 4 experimental

conditions), discourse competitor x each participant-related predictor (BPI, SES, WM,

cognitive control), visual competitor x each participant-related predictor (BPI, SES,

WM, cognitive control), BPI x SES, BPI x proficiency, WM x proficiency. Gender

was added as a covariate. Age correlated strongly with other participant-related

predictors and could therefore not be included in the models without resulting in lack

of convergence. In the following we report the optimal models.

To be consistent with the protocol in Fukumura et al. (2010) we excluded

references to the competitor. The total amount of data points expected, given the

number of participants (172) and items (20) was 3440, there were 66 no response

therefore the actual number was 3374. We excluded the following data from all

analyses: 86 items were excluded because of reference to the competitor, or

because the utterance was (partly) unintelligible. We also excluded a problematic

experimental item (N = 115) for a total of 201 items, i.e. 6% of the data.

In the first analysis, the repeated name was expected to feature as the subject

in the first sentence that participants produced to describe the second picture in the

experimental item. As in Fukumura et al. (2010) we excluded a further 155 tokens

where the target referent was indefinite or lacked a determiner, as well as 310 tokens

that were not exact repetitions of the named referent. Altogether, 19% of the data was

excluded from the first analysis. The remaining responses included a total of 1766

NPs and 942 pronouns.

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The dependent variable was the likelihood of producing a definite NP (as

opposed to a pronoun) to identify the target referent in the second picture of the

experimental items. We used logistic regression to model the probability (in terms of

logits) associated with the values of the dependent variable. NP use was predicted

by the visual presence of a competitor (z = 3.21, p

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the Simon task score and the presence of a discourse competitor (z = 2.12, p = .03).

Further, there was an interaction between language experience

(monolingual/bilingual) and language proficiency (below/above the monolingual

mean) (z = -2.15, p = .03) whereby monolingual children below the language

proficiency mean used more NPs than bilingual children below the language

proficiency mean. For children above the language proficiency mean there was no

difference as a function of language experience as shown in Figure 5.

Insert Figure 5 here

As children used NPs other than the repeated name in their story continuation,

in a second set of logistic regression analyses, we investigated the level of

informativity of the label used to identify the target referent. The dependent variable

included all the referential expressions that children used to identify a target referent

where there was evidence of an attempt at discourse integration; we therefore

excluded all bare nouns, indefinite NPs and indefinite pronouns (155 items), with

8.3% of data excluded in total. The dependent variable was binary and had two

levels: (1) underinformative expressions - third person singular pronouns and less

informative definite NPs (e.g. the man; the lady), and (2) more informative definite

NPs (repeated NPs from the preamble, semantic substitutions, e.g. the prince for the

king). Using the WM score where language proficiency and age were partialled out

did not allow the model to converge, we therefore used the raw WM score. The

optimal model shows that children were more informative in the presence of a visual

competitor (z = 2.15, p = .03), while the mention of a discourse competitor had no

significant effect (z = -1.15, p = .25). The interaction between WM and language

proficiency was a significant predictor of informativity (z = 9.59, p < .001), while

none of the other predictors made a significant contribution to the model.

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As we did earlier, we repeated this analysis including the mean monolingual

language proficiency as a threshold to investigate a potential language proficiency

disadvantage for bilingual children in the production of informative NPs. The effect

of visual competitor was significant (z = 2.14, p = 0.03), as was the effect of WM (z =

4.88, p < .001). Similarly to what we found in the first set of analyses, monolingual

children (z = 3.56, p < 0.001) and children with language proficiency above the

monolingual mean (z = 9.51, p < 0.001) produced significantly more informative NPs.

The significant interaction between language proficiency and language experience (z

= -2.18, p = 0.03) showed once again that there was no difference as a function of

language experience for children whose proficiency was above the monolingual

mean, but for those below the mean threshold monolinguals produced more

informative NPs.

Our third and final set of analyses investigated the possible causes for not

encoding the target referent with a definite NP or a pronoun (which resulted in

exclusion from the first and the second analyses). This third analysis revealed

whether children were able to integrate the discourse information provided in the

preamble – where the target was introduced with or without a competitor – and the

target in their own scene description. The dependent variable was the definiteness of

the target expression used, a proxy measure for discourse integration. Only bare

nouns were excluded (44 items), on top of the items excluded from all analyses. The

excluded items amounted to 7.3% of the data in total. In this logistic regression

analysis, the coefficients indicate the likelihood of using a definite expression, thereby

integrating the target expression with the preceding discourse without discriminating

further between more informative full NPs and less informative pronouns. Very few

items displayed lack of discourse integration: 3% in monolinguals and 4% in

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bilinguals.

The presence of a visual competitor adversely affected discourse integration (z

= -2.87, p

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discourse and visual information in a complex referential communication task.

Cognitive control skills, verbal WM, , language proficiency, language exposure and

use, and SES were investigated as predictors of the choice of discourse-appropriate

anaphoric expressions in the task.

��

With the exception of analysis 2, cognitive control – as indexed by the Simon

task score – was a significant predictor of NP use. In analysis 1 and 3 – when a

language proficiency threshold is introduced as a predictor - better cognitive control

predicted sensitivity to the presence of a discourse competitor. In analysis 3, better

cognitive control also predicted discourse integration in the absence of the additional

language proficiency threshold.

Within the context of the current experiment, the manipulation of the presence

and discourse mention of a competitor to the target referent unpredictably varied the

need to resolve a referential conflict. In the condition in which the target had no

linguistic or perceptual competition no conflict arose. However, in the remaining

three conditions the discourse and/or perceptual presence of a competitor created a

referential conflict. The resolution of this conflict required the children to both inhibit

the preferred choice of a pronoun for a recently mentioned target referent, and to use a

more informative referential expression (a NP) instead for the benefit of their

addressee. The unpredictability of an upcoming potential referential conflict

necessitated a level of monitoring that we hypothesised would correlate with their

cognitive control abilities as indexed by the performance on the Simon task.

We never found an interaction between language experience and cognitive

control in the prediction of NP use suggesting that cognitive control abilities

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conferred an advantage to both groups of children independently of bilingualism,

contrary to our initial hypothesis. This could be because the bilingual advantage for

cognitive control abilities in this group of children was modest (albeit significant, see

also De Cat et al., 2017). In our predictions we also hypothesised that whatever

bilingual advantage there might be in cognitive control might be offset by bilingual

children’s lower proficiency skills. We did find, at least in analysis 1 and 3, that the

degree of exposure and use of the home language negatively correlated with NP use

before controlling for language proficiency. In an additional set of analyses we

investigated whether keeping language proficiency constant for the monolingual and

the bilingual children might mitigate the proficiency disadvantage against the

bilinguals. Using the mean performance of the monolingual children on the language

proficiency task we split the groups above and below the monolingual mean, and we

did repeatedly found that those bilingual children that had language proficiency skills

above the monolingual mean were no different from their monolingual counterparts in

the use of informative NPs. They were however no better, as might be expected on the

assumption of a bilingual advantage in cognitive control. The reason for this lack of

bilingual advantage, once proficiency was controlled for, is likely to stem from the

heterogeneity of our bilingual group. We deliberately had very broad selection criteria

for the bilingual children in our recruitment schools so that we could include all of the

children that were classified in the UK education system as having English as an

additional language (EAL learners). This resulted in children who differed vastly in

the cumulative amount of input and output and in the range of languages spoken. As

our understanding of the bilingual cognitive advantage is progressively refined we

now know that a large number of variables, both at the level of the individual

bilingual speakers and at the level of the tasks used (Mishra et al., 2012), can

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Individual differences in the production of referential...

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